Author's Accepted Manuscript

Author's Accepted Manuscript

Author’s Accepted Manuscript Evaluating ‘‘Cash-for-Clunkers’’: Program Effects on Auto Sales and the Environment Shanjun Li, Joshua Linn, Elisheba Spiller PII: S0095-0696(12)00067-8 DOI: Reference: YJEEM1746 To appear in: Journal of Environmental Economics and Management Received date: 3 October 2011 Revised date: 9 July 2012 Accepted date: 13 July 2012 Cite this article as: Shanjun Li, Joshua Linn and Elisheba Spiller, Evaluating ‘‘Cash-forClunkers’’: Program Effects on Auto Sales and the Environment, Journal of Environmental Economics and Management, j.jeem.2012.07.004 This is a PDF file of an unedited manuscript that has been accepted for publication.

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  • Evaluating “Cash-for-Clunkers”: Program Effects on Auto Sales and the Environment1 Shanjun Li, Joshua Linn, and Elisheba Spiller Abstract “Cash-for-Clunkers” was a $3 billion program that attempted to stimulate the U.S. economy and improve the environment by encouraging consumers to retire older vehicles and purchase fuel-efficient new vehicles. We investigate the effects of this program on new vehicle sales and the environment. Using Canada as the control group in a difference-in-differences framework, we find that, of the 0.68 million transactions that occurred under the program, the program increased new vehicle sales only by about 0.37 million during July and August of 2009, implying that approximately 45 percent of the spending went to consumers who would have purchased a new vehicle anyway. Our results cannot reject the hypothesis that there is little or no gain in sales beyond 2009. The program will reduce CO2 emissions by only 9 to 28.2 million tons based on upper and lower bounds of the estimate of the program effect on sales, implying a cost per ton ranging from $92 to $288 even after accounting for reduced criteria pollutants.
  • Keywords: Stimulus, Cash-for-Clunkers, Auto Demand, CO2 emissions JEL Classifications: Q50, H23, L62
  • 1 Shanjun Li is an Assistant Professor in the Dyson School of Applied Economics and Management at Cornell University, 424 Warren Hall, Ithaca, NY, 14853, email:, phone: (607)255- 1832, fax: (607)255-9984. Joshua Linn is a Fellow at Resources for the Future (RFF), 1616 P Street NW, Washington, DC, 20036, email:, phone: (202)328-5047, fax: (202)939-3460. Elisheba Spiller is a Post-Doctoral Research Fellow at RFF, email, phone: (202)328-5147, fax: (202)939-3460. We thank Soren Anderson, Antonio Bento, Maureen Cropper, Robert Hammond, Paul Portney, Kevin Roth, and Chris Timmins for helpful comments and Jeffrey Ferris, Marissa Meir and Shun Chonabayashi for excellent research assistance. This paper supersedes RFF Discussion paper 10-39 titled Evaluating “Cash for Clunkers”: Program Effects on Auto Sales, Jobs and the Environment.
  • 1. Introduction Amid a major recession and growing concerns about the environment, many countries have adopted programs that encourage consumers to trade in their old, inefficient vehicles in exchange for more efficient ones. In the United States, the Cash-for-Clunkers program provided eligible consumers a $3,500 or $4,500 rebate when trading in an old vehicle and purchasing or leasing a new vehicle. Many other countries, such as France, the United Kingdom, and Germany, have similar programs, which generally share the same goals: to provide stimulus to the economy by increasing auto sales, and to improve the environment. The U.S. program received enormous media attention and many considered the program to be a great success; during the program’s nearly one-month run, it generated 678,359 eligible transactions and had a cost of $2.85 billion.2 But as a matter of economic theory, it is typically quite difficult to achieve multiple goals with a single policy. The large fiscal cost and public enthusiasm for these programs, and their widespread use around the world, raise the question of just how effective they are at meeting their economic and environmental goals.
  • This study estimates the composition of the fleet of vehicles that would have been sold in the absence of the program, permitting a comprehensive evaluation of the program effect on vehicle sales, the environment and economic activity. First, we examine the program effects on the quantity and composition of new vehicle sales both during the program and in the several months before and after the program. Many observers of the program were concerned that it would primarily pull demand from adjacent months, thus providing little short-term stimulus, while others believed that the program would pull demand from several years in the future (Council of Economic Advisors, 2009). Furthermore, we are interested in analyzing whether the program affected the fuel economy distribution of new vehicle sales. Because Cash for Clunkers was promoted for stimulus and environmental reasons, we focus on two types of changes in consumer behavior caused by the program: switching from purchasing low fuel-efficiency to
  • 2 Transportation Secretary LaHood declared the program to be “wildly successful” at the end of the program, while two Op-Ed articles in the Wall Street Journal on August 2nd and 3rd raised doubts about whether the program truly increased sales and stimulated the economy. They argued that the program would most likely result in the shifting of future vehicle demand to the present and could hurt the sales of other goods.
  • high fuel-efficiency vehicles, and shifting the purchase time to take advantage of the program’s incentives. Second, we evaluate the program’s cost-effectiveness in reducing gasoline consumption and carbon dioxide (CO2) emissions by comparing total gasoline consumption as well as emissions of CO2 and criteria pollutants with and without the program. There exist many federal subsidy programs aiming to reduce U.S. gasoline consumption and CO2 emissions such as tax credits for ethanol blending and income tax incentives for purchasing hybrid vehicles. Our costeffectiveness analysis permits a comparison across these different programs. The basis for these evaluations is the difference-in-differences (DID) analysis in a vehicle demand framework based on monthly sales of new vehicles by model from 2007 to 2009. The U.S. market constitutes the treatment group in the analysis. We use Canada as the control group based on two observations as well as some statistical evidence. First, Canada did not have a similar program, while nearly a dozen European countries did in 2008 and 2009. Second, the Canadian auto market is probably the most similar to the U.S. market: in both countries in recent years before the recession, about 13-14 percent of households annually purchased a new vehicle; characteristics of vehicles sold are similar; and pre-program time trends are similar. The two main identifying assumptions are that the program did not affect the Canadian vehicle market and that the differential effect of the 2008-2009 recession across the two countries’ vehicle markets did not vary over the months in 2009. We provide some evidence supporting these assumptions in Section 3.3.

The DID analysis shows that the program increased sales of vehicles that were eligible for the rebate (eligible vehicles) and lowered sales of ineligible vehicles during the program period. Furthermore, within eligible vehicles, the positive effect was larger for those with higher fuel efficiency – which yield a higher rebate under the program. The negative effect on ineligible vehicles was stronger for those that barely missed the eligibility requirement, implying that the program caused consumers to substitute from these vehicles to eligible vehicles. We find that the program reduced sales in the months before and especially after the program, and that the effect on sales weakened over time.

The empirical results thus suggest that the program shifted consumer demand from ineligible vehicles to eligible ones as well as from preand post-program periods to program periods, with the inter-temporal shift having the strongest impact.

  • With the parameter estimates from the DID analysis, we simulate vehicle sales in the counterfactual scenario of no program. We find that the program increased sales by only 0.37 million during July and August of 2009, implying that of the 0.66 million vehicles in our sample that were purchased under the program, 0.29 million would have been purchased anyway during these two months. The program effect on vehicle sales erodes further when we look at a longer time horizon: the increase in vehicle sales during June to December of 2009 was practically zero. In addition, our simulation results show that Toyota, Honda and Nissan benefited from the program disproportionally more than other firms: with a combined market share of around 38 percent before the program, they accounted for more than 50 percent of the increased sales. The U.S.-based automakers and their dealers were facing especially low revenues prior to the program, and although sales of the vehicles produced by these automakers increased by about 14 percent in July and August, over the June-December period sales increased by less than one percent. Therefore, we conclude that the program provided little economic stimulus beyond late 2009.

Based on the simulation results for vehicle sales, we estimate the differences in total gasoline consumption, CO2 emissions, and four criteria pollutant emissions (carbon monoxide, volatile organic compounds, nitrogen oxides and exhaust particulates) with and without the program. We provide the results for 12 different cases, across which parameter and behavior assumptions vary. Over the vehicles’ lifetimes, the reduction in gasoline consumption ranges from 924.5 to 2,907.3 million gallons while that in CO2 emissions ranges from 9 to 28.2 million tons. By comparison, in 2009, U.S. gasoline consumption was 141 billion gallons and CO2 emissions from passenger vehicles were 1.1 billion metric tons.

After accounting for the program’s benefit in reducing criteria pollutants, we estimate that the program’s cost of CO2 emissions reduction ranged from $92 to $288 per ton of CO2 while that of gasoline consumption reduction ranged from $0.89 to $2.80 per gallon.

Several recent studies have evaluated particular aspects of the Cash-for-Clunkers program. Knittel (2009) estimates the implied cost of the program in reducing CO2 emissions. Council of Economic Advisors (CEA 2009) and Cooper et al. (2010) analyze program impacts on vehicle sales and employment. National Highway and Traffic Safety Administration (NHTSA, 2009) also examines program effects on gasoline consumption and the environment. The major

  • difference between our analysis and the aforementioned studies lies in the fact that we use the DID approach to estimate counterfactual sales by vehicle model in the absence of the program. Knittel (2009) does not establish the counterfactual and does not examine program effects on vehicle sales. The other three studies estimate the sales effect based on heuristic rules and aggregate sales data and do not examine consumer substitution across models and over time. A recent study by Mian and Sufi (2010) is more closely related to ours in that we both establish counterfactual outcomes by exploiting variations in program exposure across different areas. Rather than using Canada as a control group, that paper uses the number of “clunkers” registered in U.S. cities prior to the program as a measure of ex-ante program exposure. Variation in this measure identifies the program effect, and the paper shows an almost identical short-term effect (July and August) to ours. They argue that by as early as March 2011, the program effect was completely reversed. Copeland and Kahn (2011) use a time-series approach to examine the program effect on sales and on production. They find a slightly larger short-term effect on vehicles sales but they also conclude that by January 2010, the cumulative effect of the program on sales was essentially zero. Neither of these papers examines environmental outcomes. Carefully analyzing the counterfactual is important for estimating the environmental benefits of the program. For example, we find a smaller cost per ton of CO2 reduction than Knittel (2009) because we account for the difference between total CO2 emissions during the remaining lifetime of the trade-in vehicles and the emissions from the new vehicles purchased to replace them, and the fact that the fleet of new vehicles purchased under the program is more fuel efficient than that without the program; Knittel (2009) only considers the first effect. Failing to analyze the counterfactual fleet without the program can thus underestimate the program’s environmental benefit.

2. Background and Data In this section, we first discuss the background of the Cash-for-Clunkers program, including the timeline and eligibility rules. Next, we present the data that are used in the empirical analysis. 2.1 Program Description

  • As Figure 1 shows, the Consumer Assistance to Recycle and Save (CARS) Act was passed by the House of Representatives on June 9th , 2009 and by the Senate on June 18th , and was signed into law by the President on June 24th . The law established the Cash-for-Clunkers program, a temporary program granting subsidies to individuals who trade in their older, fuel inefficient, vehicle to purchase a new and more efficient vehicle. The traded-in vehicle would then be dismantled in order to ensure that it does not return to the road. The program was officially launched on July 27th , 2009 and terminated ahead of schedule on August 25th , 2009. It generated 678,359 eligible transactions at a cost of $2.85 billion.3 Originally, the program was planned as a $1 billion program with an end date of November 1st , 2009.
  • Figure 1: Timeline of the Cash-for-Clunkers Program June 9 June 24 July 27 August 25 “C-f-C” President CARS program CARS program approved signed CARS officially launched ended by House by NHTSA June 18 July 24 November 1 Bill approved Final rules Projected by Senate issued end date
  • The Cash-for-Clunkers program was intended to reduce the number of old and less fuel efficient vehicles (i.e. clunkers) on the roads as well as shift demand towards more fuel efficient new vehicles. The program outlined four requirements that the trade-in would have to meet in order to be eligible, as shown in Table 1A. These requirements varied according to the size and class of the vehicle. The first three requirements ensured that the traded-in vehicle would otherwise be on the road had it not been for the program: the trade-in vehicle must be drivable; it must have been continually insured and registered by the same owner for the past year; and it must be less than 25 years old. The fourth rule ensured that the vehicle is in fact a “clunker”: it
  • 3 Statistics are from press releases at Pre-Program Period Program Period PostProgram Period
  • must have a combined fuel efficiency of 18 mpg or less (the latter two requirements were different for category 3 trucks).4 Table 1B shows the minimum MPG a new vehicle needed to qualify. The MPG requirement was 22 for passenger automobiles, 18 for category 1 trucks, and 15 for category 2 trucks. Category 3 trucks, on the other hand, had no minimum fuel efficiency requirement, but they could only be traded in for category 3 trucks. Finally, the manufacturer’s suggested retail price (MSRP) of the new vehicle could not exceed $45,000. Table 1B shows that the stringency of the MPG requirement was greatest for passenger cars and decreased across the truck categories. For example, a new passenger car must have an MPG improvement of at least 4 over the trade-in vehicle in order to qualify for the $3,500 rebate while a 10 MPG improvement is needed for the $4,500 rebate. For a new vehicle in category 1, the requirements on the MPG improvement are 2 and 5 for the two rebate levels. The requirements become even less stringent for category 2 and 3 vehicles.

2.2 Data Description We collect data on monthly vehicle sales for all models in the United States and Canada from 2007 to 2009 from Automotive News. We combine these data with vehicle MPG data from the Environmental Protection Agency’s fuel economy database as well as vehicle prices and other characteristics from Wards’ Automotive Yearbook. Our data include 16,776 observations of monthly vehicle sales. We define a model as a country-vintage-nameplate (e.g., a 2007 Toyota Camry in the United States) and we have 1,436 models in the data. Almost all models sold in Canada are available in the United States.

  • Table 2 provides summary statistics of the data set. Based on the eligibility rules, 1,008 of the 1,436 vehicle models meet the requirement and could be eligible for the rebate during the program (henceforth, eligible vehicles). Among the 16,776 observations, about 70 percent of sales in both countries are for eligible vehicles. As shown in the table, the eligible vehicles have much higher sales than ineligible ones. Although average sales per model in the United States are
  • 4 Category 1 trucks are “non-passenger automobiles” including SUVs, medium-duty passenger vehicles, pickup trucks, minivans and cargo vans. Category 2 trucks are large vans or large pickup trucks whose wheelbase exceeds 115 inches for pickups and 124 for vans. Category 3 trucks include very large pickup trucks and cargo vans.
  • much higher than in Canada, the number of new vehicles sold per household is 13-14 percent in both countries. On average, the eligible vehicles are cheaper and, by definition, more fuelefficient than the ineligible ones. The average prices (sales-weighted) are very similar in the two countries across both categories. Because the share of light trucks in total sales is larger in Canada than the United States, the average fuel efficiency of vehicles in Canada in both categories is lower than in the United States.

To examine the effectiveness of the program on energy consumption and the environment, we use the public database for the Cash-for-Clunkers program from

The data set provides (dealer-reported) information on the trade-in and new vehicles for each transaction during the program. There are 678,539 transactions in the data set. We remove transactions that are subject to reporting error (e.g., reported MPG that does not meet the eligibility criteria). In addition, we delete 2,278 category 3 vehicles and 6,169 leased vehicles in order to be consistent with our demand analysis of new vehicles. After removing 18,959 records, there are 659,400 observations of trade-in and new vehicles under the program.

Table 3 shows the summary statistics on trade-in and new vehicles. This table demonstrates that consumers were trading in more light trucks than cars, and that these trucks were newer than the cars. Our final sample has an average rebate amount of $4,214 and a total payment of $2.78 billion (out of $2.85 billion for all transactions in the full data set). 3. Empirical Strategy In this section, we first discuss the channels through which the program could affect vehicle sales. We then describe our empirical model.

3.1 Potential Program Effects In our analysis, we assume that the program did not affect vehicles sales prior to June 2009.

Although some consumers may have known about the bill before the House passed it on June 9th , we expect that the uncertainty surrounding the eligibility requirements as well as the bill’s final passage would greatly limit its effect before June 9th . In fact, our estimation results show that there is no significant effect on sales even in June. The program period is defined from July 27th

  • to August 25th . Although the program retrospectively recognized qualified sales from July 1st until the official start date, the total number of these pre-program sales was only 30,317, which is less than the average daily sales during the first week of the program. Because an automobile is a durable good, the program could affect vehicle sales before, during, or after the program period. During the program period, some consumers who would have purchased an ineligible model or chosen not to purchase a new vehicle may choose to purchase an eligible model instead, as depicted in Figure 2. In addition, the program could result in consumers changing the purchase time in order to coincide with the program period (i.e., intertemporal substitution). In the absence of the program, these consumers could have purchased an eligible or an ineligible vehicle in other periods. Both channels would increase total vehicle sales and likely improve fleet fuel-efficiency. To a large extent, the design of the program in achieving the stimulus purpose was to pull demand forward from a sufficiently distant future when the economy was expected to be stronger. Thus, the time horizon over which the intertemporal substitution occurs is crucially important to the stimulus purpose but not so for the environmental purpose. The graph below illustrates the different substitution channels. Figure 2: Diagram of Program Effects Choices \ Timing Pre-program 06/01-07/26 Program 07/27-08/25 Post-program 08/26- Ineligible Vehicle Eligible Vehicle No Purchase The degree of these substitutions could vary over product space as well as over time for several reasons. First, there could be a stronger substitution to eligible vehicles from vehicles that barely miss the MPG requirement, compared to the substitution from vehicles that have much lower fuel efficiency. This is due to the fact that higher fuel-efficiency vehicles tend to
  • compromise on certain amenities such as horsepower and engine size, and thus a consumer would face a smaller trade off in amenities by only marginally increasing fuel efficiency. In addition, because high MPG vehicles could be eligible for a higher rebate ($4,500 versus $3,500) the program could have a stronger effect on the vehicles eligible for the higher rebate. Second, the substitution could exhibit heterogeneity over time. Intuitively, the intertemporal substitution should be stronger right before or after the program than farther away from the program. Moreover, because the length of the program is not fixed and runs out when the designated amount of stimulus money is used up, the program could have a stronger stimulus effect at the beginning of the program period. In fact, the initial one billion dollars were used up within a week while the additional two billion dollars lasted for three weeks. Thus, we explicitly model and measure these substitution patterns in our estimation.

3.2 Empirical Model We implement the DID method in a regression framework where the Canadian auto market is used as the control group for the U.S. market. Our DID regression estimates how the program affected vehicle sales before, during, and after the program period on a monthly basis given the vehicle’s eligibility and other characteristics. The causal interpretation hinges on the identifying assumption that (unobserved) demand and supply shocks at the time of the program are the same across the two countries. Section 3.3 presents analysis suggesting that Canada is a valid control group to estimate underlying trends that are not affected by the program but that do affect vehicle sales (such as economic shocks that occur at the same time as the program).

The regression model is based on monthly sales of new vehicles by vehicle model. Let c index country (United States or Canada), t index year, m index month, and j index vehicle nameplate (e.g., Ford Focus). We define a vehicle model as a country-year-nameplate (e.g., a 2009 Ford Focus in the United States) and use ctj as the index. By including interactions of month dummies with eligibility in a regression framework, the program can have different effects across months and eligibility status. This allows us to identify both the intertemporal and cross-model substitution patterns discussed in the previous section.

  • We define ctj E as the eligibility dummy, equal to one for any vehicle in either country that meets the program requirement (irrespective of whether the program is in effect) and zero otherwise. ctj I is a dummy variable for ineligible vehicles and is equal to one for any vehicle in either country that does not meet the program requirement. ctm P is a dummy variable equal to one for months when the program may have had an effect (e.g., June to December of 2009) in the United States and zero otherwise. The interaction of the program dummy with eligibility dummies indicates models that are in the program and are eligible for the rebate. Equation (1) allows us to disentangle monthly program effect on sales for eligible and ineligible vehicles.
  • E I ctmj ctj ctm tm ctj ctm tm log ctmj ctj ctj cm ctj cm ctj tm ctj tm ctmj 1) where ctmj q is the sales of vehicle model j.5 The first term on the right side captures the program effect on eligible vehicles during the relevant months while the second term captures that on ineligible vehicles. Instead of estimating the program effects for pre-program, during-program, and post-program periods, we estimate the effects month by month using this flexible specification for two reasons: (1) the program period does not coincide with a full month and our data are at the monthly level; and (2) the nature of intertemporal substitution would imply a diminishing impact farther away from (either before or after) the program period. The first two terms capture the program effect on vehicle sales in the United States and these two terms are zero for the observations in Canada. However, interpreting these coefficients as causal program effects hinges on the assumption that Canada is a valid control group. The other variables in the equation help identify the impact of the program on sales by controlling for observed and unobserved country and vehicle attributes.
  • Because the program affected demand for eligible vehicles in proportion to their fuel efficiency, we must control for the effect on sales of fuel costs, which also depends on fuel efficiency.
  • 5 For all the regressions presented in the paper, we also estimate a multinomial logit model in the linear form (Berry 1994) where we assume that consumers have a total of J vehicle models plus an outside good indexed by 0 (i.e., not purchasing a new vehicle) to choose from in a given month. The dependent variable is 0 log( ) log( ) ctmj ctm S S
  • with ctmj S and 0 ctm S being the market shares of model j and the outside good that captures the decision of not purchasing a new vehicle. The market size is the number of households in the two countries. The results are very close to the results from the linear models shown in Section 4.
  • Variables in ctmj x include dollars per mile (gasoline price/MPG), which is proportional to the lifetime fuel costs of the vehicle assuming the price of gasoline follows a random walk. Several recent empirical studies have documented a negative relationship between fuel costs and vehicle sales (e.g., Busse et al. 2009, Li et al. 2009, and Klier and Linn 2010). We allow the coefficient on dollars per mile to be different in these two countries. Since we control for model (countryyear-nameplate) fixed effects, vehicle MPG itself is subsumed in these fixed effects. ctj [ denotes model (i.e., country-year-nameplate) fixed effects, which control for monthinvariant observed and unobserved vehicle attributes (such as horsepower, weight, and product quality), as well as month-invariant demand shocks at the model level. E cm K and I cm K are countrymonth fixed effects to capture country-specific seasonality for eligible and ineligible vehicles (such as the December holiday effect , E ctj cm [ K and I cm K are all country-specific fixed effects, controlling for country-specific demand and supply shocks that affect the level of vehicle sales (these would be equivalent to household or firm dummies in a canonical DID example). Because the fixed effects vary by nameplate-year-country, we allow for the possibility that the recession or other shocks affected the U.S. and Canadian vehicle markets differently each year. E tm G and I tm G are year-month fixed effects for eligible and ineligible vehicles (these would be equivalent to time dummies in a canonical DID example) common across countries. Because these fixed effects are used to capture demand shocks for the two groups of vehicles that are common in the two automobile markets, they give rise to the control group interpretation for the Canadian market.6 Finally, ctmj H is the random demand shock. Although equation (1) provides a starting point for our analysis, it does not allow heterogeneous program effects across vehicles within the same eligibility category. Heterogeneous effects could exist among both eligible and ineligible vehicles for the reasons noted in the previous section. First, since there are two rebate levels ($3,500 and $4,500) and the size of the rebate depends on the difference between the MPG of the new vehicle and that of the trade-in vehicle, consumers may substitute towards eligible vehicles with higher MPGs as these vehicles are more likely to provide them with a $4,500 rebate. Second, the program effect on ineligible vehicles could be correlated with fuel efficiency as well: consumers are more likely to switch from barely ineligible vehicles to eligible vehicles, rather than substitute away from vehicles much farther from the eligibility cut-off. Due to the trade-offs between vehicle size/horsepower and fuel efficiency, consumers likely suffer a smaller sacrifice in vehicle size or horsepower by switching from barely ineligible vehicles to eligible ones, rather than from
  • 6 Because not all models are available in both countries, we cannot use year-month-model fixed effects.
  • vehicles that are far below the MPG requirements. To capture the heterogeneous effect, we estimate equation (2) as our main specification.
  • E E ctmj ctj ctm tm ctj ctm ctj tm log GPM ctj ctm tm ctj ctm ctj tm I P I P GPM ctmj ctj ctj cm ctj cm ctj tm ctj tm ctmj 2) GPM is gallons per mile and * 1/ 1/ j j GPM MPG MPG
  • C , where * MPG is the MPG requirement for rebate eligibility, which varies across vehicle categories as discussed in Section 2.1. Thus,
  • ctj GPM measures how far a vehicle’s fuel efficiency is from the eligibility requirement. The farther away a vehicle’s fuel efficiency is from the requirement (for either an eligible or ineligible vehicle), the larger the variable is. The second and fourth terms on the right side of the equation capture the heterogeneous program effect for vehicles within the same eligibility category. As the results show below, although equations (1) and (2) provide similar estimation results for the program effect on vehicle sales, equation (2) leads to a much larger effect on average vehicle fuel economy (and hence a larger environmental benefit and energy savings).

3.3 Canada as the Control Group In this section, we provide qualitative and quantitative support for using Canada as the control group. First, Canada did not have a similar program, whereas many European countries including Germany, France, Italy and Spain did in 2008 and 2009. Although Canada has a Retire Your Ride Program that started in January 2009, the program is not comparable to the Cash-forClunkers program for at least three reasons. First, the program provides only CA$300 worth of credit for eligible participants (owners of pre-1996 model-year vehicles that are in running condition), compared to $3,500 or $4,500 offered in the United States.

Second, the goal of the Canadian program is to improve air quality by encouraging people to use environmental-friendly transportation, so the program is not tied to new vehicle purchases. Depending on the province, the credit can be a public transit pass, a membership to a car-sharing program, cash, or a rebate on the purchase of a 2004 or newer vehicle. Third, the program only retired about 60,000 vehicles during the first 15 months. Therefore, its effect on new vehicle sales (about 1.6 million annually) should be negligible.

  • The second justification for using the Canadian auto market as the control group is that it is probably the most similar to the U.S. market. About 13-14 percent of households purchased a new vehicle in recent years before the economic downturn in both countries. Table 2 also shows that the vehicles sold have similar characteristics, although the U.S. market has a larger set of models. Figure 3 depicts monthly sales in logarithm of all, eligible, and ineligible new vehicles in the two countries from 2007 to 2009. By and large, the two series track each other well. A noticeable difference is that sales in Canada seem to have stronger seasonality (e.g., a larger hump during March-May each year), suggesting the importance of controlling for countryspecific seasonality in our analysis.

Our empirical models given in equations (1) and (2) control for unobservables in several dimensions by including model fixed effects ctj [ , common year-month fixed effects tm G , and country-specific seasonality cm K . Nevertheless, as we discussed above, the unbiasedness of the coefficient estimates hinges on the identifying assumption that the time trends in demand and supply are the same in the two countries. Otherwise, we risk interpreting preexisting differences in time trends as the effect of the program.

  • The economic downturn that started in the second half of 2008 raises a particular concern that the demand and supply trends were not similar in the two markets. The recession in the United States was driven by the housing market crisis; the mortgage default rate increased dramatically and housing prices fell sharply at the onset of the crisis. By comparison, housing prices in Canada continued to increase until late 2008. In addition, the credit market in Canada was not impaired and did not experience the same “credit crunch” as the United States. As a result, the downturn in Canada was milder and the auto market in Canada did not contract as much as in the United States. 7,8 It is important to note that because equation (2) includes nameplate-country-year fixed effects, our model allows for the possibility that the recession differentially affected the U.S. and Canadian markets. For example, the estimated program effects would not be biased if the auto
  • 7 We examine the trends of durable goods sales across both countries during this time and find similar pre-trend declines in spending across electronics, appliances, and new vehicles.
  • 8 Although GDP growth, employment, and household spending slowed in Canada, the decrease in total new vehicle sales in the second half of 2008 was less severe in Canada: sales dropped 1.1% in Canada in 2008 (against a 1.5% increase in 2007) while they dropped 18% in the United States (against a 3% drop in 2007).
  • market contracted in 2008 to different degrees in the U.S. and Canada. Nevertheless, we must maintain the assumption that the differences across countries in the effects of the recession do not vary over the year, which raises two concerns. First, we control for seasonality by including country-month fixed effects, but the estimated fixed effects would be biased if, for example, the downturn in August 2008 was more severe than in other months in 2008; in turn this would bias the estimated program effects. To address this concern, we drop the data from June to December of 2008 as an alternative to the estimation using the full data set. If the downturn were causing significant bias, we would expect to obtain different results by omitting these observations. As we show below, we obtain qualitatively similar results from these two estimations. The second concern about the recession is that the effects of the recession immediately before, during, or after the program may have been different than the effects at other times during 2009. In the difference-in-difference framework, we maintain the identifying assumption that the relative effects of the recession on the United States and Canada were similar throughout 2009. If the recession had a larger negative effect on the U.S. market during the months prior to the program than during the program, the estimated program effects would be biased away from zero. This identifying assumption cannot be directly tested, but we can take advantage of the data before the program period to examine differences in pre-existing trends. Similarity before the program would support the assumption that the trends are the same during and afterwards. To that end, we estimate equation (2) without the first four terms on the right hand side using data before June 2009 (i.e., before the program affected the market). Figure 4 plots the aggregate monthly sales after removing the effects from observed variables ( ) ctmj x , time trends tm G , and seasonality cm K . The three panels show that underlying vehicles sales (i.e., residuals) track each other quite well in the two countries before the program.

To gauge the importance of using the Canadian market as the control group, we estimate a model without the control group and present the results in the online appendix (Appendix Tables 2 and 3) posted at the journal’s online repository of supplemental material, which can be accessed via We find much larger program effects on vehicle sales from this analysis than the DID analysis in both the short and long run. In addition, we do not find that the positive sales effect erodes over time, suggesting a lack of evidence of intertemporal substitution. This demonstrates the importance of having a valid control group; ignoring the underlying trend biases the results and causes the program to look much more successful than it truly was.

Thus, in the main text, we focus on the DID analysis as our main specification.

  • 4. Estimation Results We first present parameter estimates for equations (1) and (2). We then discuss program effects on vehicle sales and fuel economy implied by these parameter estimates. 4.1 Difference-in-Differences Results Table 4 reports parameter estimates and standard errors for three regressions. The first regression is equation (1) while the second one is equation (2), both using the full sample. The third estimation is for equation (2) based on the sample without the second half of 2008 (shorter sample). We only report the coefficient estimates associated with program effects (June to December of 2009) for the two groups of vehicles, noting that the full set of control variables described after equation (1) is included in the regressions; estimates of the other coefficients have the expected signs and are available upon request.9 In the second and third regressions, we include the interaction of the vehicle eligible dummy and |
  • GPM| to allow for heterogeneous effects across vehicles.10 For example, in the top panel, the first row shows the effect of the program on sales of eligible vehicles in June, i.e., before the program begins. The second row shows whether the effect is larger for vehicles that are further from the MPG requirement. Subsequent rows show analogous coefficients for other months, and the bottom panel reports coefficient estimates for ineligible vehicles. Throughout the paper, standard errors are constructed using block bootstrap and are robust to heteroskedasticity and serial correlation within a vehicle model (country-year-nameplate). We also estimate standard errors with two alternative block definitions: country-nameplate and nameplate. The standard errors are slightly smaller under both alternative definitions. As a conservative measure, we present the standard errors using a vehicle model as a block in the main text. The alternative standard errors for the simulations of program effects are presented in the online appendix (Appendix Table 4). Overall, the parameter estimates have the expected signs. The directions of the program effect on sales suggested by the parameter estimates are similar across all three estimations, and
  • For the second regression, the coefficient estimates on dollars per mile are -9.877 (1.249) for Canada, and -10.075 (1.254) for the United States. The estimates are negative as well in the other two regressions. 10 The mean of |
  • GPM| for eligible vehicles in 2009 in the United States is 0.67 with a range from 0 to 2.61. The mean of |
  • GPM| for ineligible vehicles is 0.67 with a range from 0.21 to 1.70.
  • we focus on the full-sample results. The fourth column in the top panel in Table 4 shows the parameter estimates using the full sample. The two coefficient estimates for June suggest that the program reduced sales of eligible vehicles but the reduction is smaller for high MPG vehicles, both without statistical significance. The two coefficient estimates for July capture the combined effects from the pre-program period (July 1st -26th ) and the program period (27th -31st ). We would expect a decrease in sales during the pre-program period and an increase during the program period. Therefore, the combined effect could be positive or negative. The coefficient estimates using the full sample suggests that the program reduced the sales of eligible vehicles with low MPG while it increased the sales of those with high MPG. Similarly, the coefficients for August capture the combined effect during the program (August 1st -25th ) and post-program (August 26th - 31th ). The coefficient estimates imply that the combined effect on eligible vehicles was positive and that the increase in sales was larger for eligible vehicles with high MPG. These results imply that the positive program effect outweighed the negative intertemporal substitution effect in both July and August.
  • The coefficient estimates for September suggest that the program reduced sales of eligible vehicles and that the decrease in sales was larger for eligible vehicles with high MPGs, consistent with consumers moving purchases forward to take advantage of the program. The parameter estimates for October and November suggest a negative effect on sales but the estimates are not statistically significant. For ineligible vehicles, the parameter estimates suggest a negative effect from July to December and a larger effect for vehicles that miss the MPG requirement by a smaller margin (e.g., a smaller |
  • GPM|). This is consistent with the fact that when consumers switch from these vehicles to eligible vehicles, they do not need to make a large sacrifice in other vehicle attributes such as horsepower and size, as discussed in Section 3. The third column shows that the results are qualitatively similar for the short sample, both for eligible and ineligible vehicles. It is important to point out that our empirical model assumes that there are no interactions between the two markets (e.g., the Cash-for-Clunkers program does not affect the Canadian market). Sales may be correlated across countries for a variety of reasons, but a particular concern for the empirical strategy would be if demand in the U.S. affects the availability or prices of vehicles in Canada. For example, during the program the greater U.S. demand for
  • eligible vehicles could cause manufacturers to divert to the U.S. market eligible vehicles that would otherwise have been supplied to Canada.11 This would decrease sales of eligible vehicles in Canada, and potentially bias the estimated program effects away from zero. In that sense we consider the reported estimates to be an upper bound, which strengthens the main conclusion that the program had a very small effect on total sales.12 4.2 Program Effect on New Vehicle Sales and Fuel Efficiency Based on the parameter estimates from Table 4, we simulate new vehicle sales under the counterfactual scenario without the Cash-for-Clunkers program. The two plots in Figure 5 show sales effects for eligible and ineligible vehicles from June to December of 2009 for the full sample based on parameter estimates from equation (2). Dashed curves represent the 90 percent confidence intervals estimated by bootstrap. The point estimates show the differences between observed and simulated sales. The corresponding plots in Figure 6 are based on parameter estimates using the short pre-program sample.
  • The results in both figures demonstrate the two channels through which the program affects vehicles sales (as discussed in Section 3.1). First, the sales of eligible vehicles increased in July and August but decreased in adjacent months, implying that some consumers shifted their purchase timing. Second, the program had a strong positive effect for eligible vehicles in August but a negative effect for ineligible vehicles from July to December, especially in August, suggesting that some consumers switched from ineligible vehicles to eligible vehicles. The effect on sales in June was negative but not statistically different from zero in both estimations, supporting our modeling assumption that the program effect before June was negligible. Because the program was implemented from July 27th -August 25th , the effect on total sales in July and August captures the (positive) effect during the program period and the (negative) effect due to intertemporal substitution just before or after the program. The net effects are both positive in July and August, although the effect in July is not statistically
  • 11 The abrupt start of the program, the short program-period, and the large vehicle inventories before the program started make this less likely to have occurred.

12 Another interaction between the two countries can occur if manufacturers made strategic pricing decisions across both countries, thus the boom in demand due to the program in the United States could potentially affect supply in Canada. However, the short timing of the program and the relatively small amount of extra vehicles sold during the period most likely mitigates this effect.

  • significant in the second estimation. The sales effects are all negative in September to November from both figures, particularly in the second estimation. Figure 7 shows the cumulative effects over different time horizons. The left-most point shows the cumulative effect during July-August. The points to the right show that the positive effects eroded over time. The top plot (based on the full sample) shows that the net effect is not statistically different from zero by the end of October. The bottom plot (based on the short preprogram sample) shows the same result by the end of September. Both plots show that the program likely had a short-lived effect on total vehicle sales. Panel 1 of Table 5 reports monthly observed and simulated sales of new vehicles from June to December of 2009. Column (1) gives the observed sales while columns (2) and (3) provide the estimated sales effects and standard errors based on the parameter estimates from equation (1) using the full sample. Columns (4) to (5) provide results based on the parameter estimates from equation (2) using the full sample. Columns (6) and (7) are results using the short pre-program sample.
  • The estimated sales effects are similar across all three regressions. The cumulative effect on sales during July and August is estimated to range from 345,000 units to 405,000. Specification two provides an estimate of 370,000, in the middle of the range. This suggests that out of the 660,000 program participants, about 290,000 would have purchased a new vehicle during July and August even without the program. This underscores that one cannot take the number of vehicles sold through the program as the net program effect on vehicle sales. In addition, the estimate suggests that about 45 percent of the total spending ($1.4 billion) went to consumers who would have purchased a new vehicle anyway. Looking at a longer horizon, neither of the estimates suggests a net gain in sales during the period from June to December. Our estimate of the short-term effect on sales of about 360,000 is essentially identical to that of Main and Sufi (2010), despite the fact that different control groups are used. The point estimate is smaller than the 450,000 units from Copeland and Kahn (2011), but their estimate is within the 90 percent confidence interval of ours. In addition, all three studies broadly conclude that the program effect on sales is short-lived, with ours suggesting an even shorter effect.13
  • 13 Copeland and Kahn (2011) argue that Canada had a milder downturn than the United States and as a result the rebound in the second half of 2009 could be milder as well. If our model was not able to address
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